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British Journal of Psychology (2011), 102, 443–463
C 2011 The British Psychological Society
The
British
Psychological
Society
www.wileyonlinelibrary.com
Conspiracist ideation in Britain and Austria:
Evidence of a monological belief system and
associations between individual psychological
differences and real-world and fictitious
conspiracy theories
Viren Swami1,2 ∗ , Rebecca Coles1 , Stefan Stieger3 ,
Jakob Pietschnig3 , Adrian Furnham4 , Sherry Rehim1
and Martin Voracek3
1
Department of Psychology, University of Westminster, London, UK
Department of Psychology, HELP University College, Kuala Lumpur, Malaysia
3
Department of Basic Psychological Research, School of Psychology, University of
Vienna, Vienna, Austria
4
Department of Clinical, Educational, and Health Psychology, University College
London, London, UK
2
Despite evidence of widespread belief in conspiracy theories, there remains a dearth of
research on the individual difference correlates of conspiracist ideation. In two studies,
we sought to overcome this limitation by examining correlations between conspiracist
ideation and a range of individual psychological factors. In Study 1, 817 Britons indicated
their agreement with conspiracist ideation concerning the July 7, 2005 (7/7), London
bombings, and completed a battery of individual difference scales. Results showed that
stronger belief in 7/7 conspiracy theories was predicted by stronger belief in other realworld conspiracy theories, greater exposure to conspiracist ideation, higher political
cynicism, greater support for democratic principles, more negative attitudes to authority,
lower self-esteem, and lower Agreeableness. In Study 2, 281 Austrians indicated their
agreement with an entirely fictitious conspiracy theory and completed a battery of
individual difference measures not examined in Study 1. Results showed that belief in the
entirely fictitious conspiracy theory was significantly associated with stronger belief in
other real-world conspiracy theories, stronger paranormal beliefs, and lower crystallized
intelligence. These results are discussed in terms of the potential of identifying individual
difference constellations among conspiracy theorists.
∗ Correspondence should be addressed to Dr. Viren Swami, Department of Psychology, University of Westminster, 309 Regent
Street, London W1B 2UW, UK (e-mail: [email protected]).
DOI:10.1111/j.2044-8295.2010.02004.x
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Viren Swami et al.
Although there remains some debate as to its precise definition (see Bale, 2007; Swami
& Coles, 2010; Sunstein & Vermeule, 2009), conspiracist ideation is usually described
as a belief in the existence of a ‘vast, insidious, preternaturally effective international
conspiratorial network designed to perpetrate acts of the most fiendish character’
(Hofstadter, 1966, p. 14). Defined in this way, conspiracy theories appear to be relatively
widespread: recent surveys have shown, for example, that up to a quarter of respondents
disbelieve official accounts of the September 11, 2001, and July 7, 2005, terrorist attacks
in the United States and Britain, respectively (e.g., Hargrove & Stempel, 2006; Soni, 2007).
Given such figures, the lack of robust empirical research on the topic of conspiracy
theories comes as a surprise (Swami & Coles, 2010). Early work in the area was heavily
influenced by Hofstadter’s (1966) essay on the ‘paranoid style’, in which the delusional
nature of conspiracy theories was emphasized. Hofstadter (1966) further argued that
conspiracy theories offered a voice to those who felt powerless or disadvantaged,
particularly in the face of crisis and when mainstream accounts contained erroneous
or unreliable information (see also Leman, 2007; Miller, 2002; Whitson & Galinsky,
2008). In this view, a conspiracy theory ‘provides a convenient alternative to living with
uncertainty’ (Zarefsky, 1984, p. 72), and recent work has conceptualized conspiracist
ideation as a logical response that arises because individuals have little access to accurate
information (Hardin, 2002; Sunstein & Vermeule, 2009).
Building on Hofstadter’s (1966) early work implicating powerlessness in the aetiology
of conspiracy theories, other scholars have suggested that such ideation arises because of
an inability to attain specific goals (Edelman, 1985; Inglehart, 1987) or because they afford
a means of self-esteem maintenance (Robins & Post, 1997), coping with persecution
(Combs, Penn, & Fenigstein, 2002), reasserting individualism (Davis, 1969; Melley,
2000), or expressing negative feelings (Ungerleider & Wellisch, 1979). Where Hofstadter
(1966) argued that conspiracy theories stemmed from ‘distorted’ thinking, these recent
conceptualizations generally concur that conspiracy theories are a rational attempt to
understand complex phenomena and deal with associated feelings of powerlessness (see
Sanders & West, 2003).
Beginning in the early 1990s, psychologists began to examine the socio-cognitive
bases of conspiracy theories (e.g., Butler, Koopman, & Zimbardo, 1995; Douglas &
Sutton, 2008). Clarke (2002), for example, discussed conspiracy theories in the context
of the fundamental attribution bias: because of the general tendency to overestimate
the importance of dispositional factors and underestimate situational factors, conspiracy
theorists are more likely to blame conspiratorial agents even when there are adequate
situational explanations of an event. This bias may be heightened when individuals
experience intense emotions triggered by catastrophic events, which in turn aid the
dissemination of conspiracy theories that provide a justification for those affective states
(Sunstein & Vermeule, 2009).
McHoskey (1995) has likewise discussed conspiracy theories in the context of biased
assimilation of information and attitude polarisation. Specifically, McHoskey (1995)
showed that individuals tended to uncritically accept evidence that was supportive of
their own argument, while discrediting contrary evidence. On the other hand, when
participants were presented with mixed evidence, they tended to polarize in their
attitudes, with greater acceptance of their original viewpoint rather than a reversal
of their beliefs. More recently, Leman and Cinnirella (2007) showed that conspiracy
theorists judged fictitious accounts of an assassination more plausible if it was consistent
with their beliefs, a finding that was explained in terms of a ‘confirmation bias’.
The latter finding links well with a small body of work that has focused specifically
on the individual difference antecedents of conspiracist ideation. In a seminal study,
Conspiracist ideation
445
Goertzel (1994) argued that conspiracy beliefs form part of a ‘monological belief system’
in which a conspiratorial idea serves as evidence for other conspiracist ideation. Thus,
for example, recent work has shown that respondents who more strongly endorsed
conspiracy theories about the September 11, 2001 (9/11) terrorist attacks were more
likely to believe in other, unrelated conspiracy theories (Swami, Chamorro-Premuzic, &
Furnham, 2010). In Goertzel’s (1994) view, monological belief systems afford believers
accessible explanations for new phenomena that are difficult to explain or that threaten
existing belief systems.
Other work on individual differences and conspiracy theories has generally supported
sociological work on the topic (for a review, see Swami & Coles, 2010). For instance,
some work has shown significant associations between conspiracist ideation and greater
anomie, distrust in authority, political cynicism, powerlessness, and lower self-esteem
(Abalakina-Paap, Stephan, Craig, & Gregory, 1999; Goertzel, 1994; Swami et al., 2010).
Related work has also shown that conspiracist beliefs are associated with a higher
authoritarian tendency (Abalakina-Paap et al., 1999; McHoskey, 1995), which has been
explained as a function of the tendency among conspiracy theorists to blame outgroups
for problems or crises experienced by the ingroup.
In perhaps the most robust empirical investigation to date, Swami et al. (2010)
examined the individual difference correlates of 9/11 conspiracy theories among
a representative sample of British respondents. They reported that belief in 9/11
conspiracy theories was strongly associated with belief in other conspiracy theories
and exposure (either directly or vicariously) to 9/11 conspiracist ideation. In addition,
they also reported significant associations between 9/11 conspiracist ideation and more
negative attitudes towards authority, higher political cynicism, and greater support for
democratic principles (SDP). Finally, Swami et al. (2010) also showed that there were
significant associations between conspiracist ideation and the Big Five personality factors
of Agreeableness (a negative association, explained as a function of the relationship
between disagreeableness and suspicion and antagonism towards others) and Openness
to Experience (a positive association, explained as a function of the relationship
between openness and intellectual curiosity, active imagination, and proclivity for novel
ideas).
Although the study by Swami et al. (2010) highlighted a number of significant antecedents of conspiracist ideation, the authors also pinpointed two important limitations
to their study, which the present work sought to overcome. First, Swami et al. (2010)
cautioned that they employed a British sample to study beliefs about an event in the
United States, raising questions about the generalizability of their findings. In Study 1,
therefore, we replicated and extended the study by Swami et al. (2010) by examining
individual difference correlates of the July 7, 2005, London bombings among a British
sample. Second, Swami et al. (2010) suggested that theirs was a preliminary study that
did not exhaust the potential list of individual difference factors that may be associated
with conspiracist ideation. In Study 2, therefore, we examined the association between
an entirely fictitious conspiracy theory specific and a range of previously unexamined
individual difference factors.
STUDY 1: 7/7 CONSPIRACIST IDEATION
The London bombings on July 7, 2005 (7/7) were a series of coordinated suicide attacks
on London’s public transport system, carried out by four British Muslim men who
appeared to be motivated by Britain’s involvement in the wars in Afghanistan and Iraq.
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Viren Swami et al.
In the absence of a public inquiry into the bombings, however, a number of conspiracy
theories have postulated alternative explanations of the events. Although there are a
number of different variants of such conspiracy theories, most share the basic claim that
the British public have not been told the whole truth about the 7/7 bombings and that
the British government either perpetrated or allowed the attacks to happen in order to
increase support for extending the wars in Afghanistan and Iraq and clamping down on
civil liberties domestically (e.g., Obachike, 2007).
More specific claims include the suggestion that the bombers were patsies for British
intelligence (based on inaccuracies about timings of trains between the bombers’ point
of departure and destination), claims of explosions under (rather than in) train carriages,
and allegations of doctoring or faking of photographs of the bombers. Indeed, such claims
appear to have relatively widespread mileage, as documented in the British Broadcasting
Company’s documentary series The Conspiracy Files. Moreover, a survey of 500 British
Muslims reported that more than half of the respondents did not believe that the whole
truth about the 7/7 events had been disclosed and that British intelligence services
doctored evidence to convict terrorist suspects (Soni, 2007).
Clearly, then, there are important reasons for examining the individual difference
correlates of beliefs in 7/7 conspiracy theories (see also Swami & Coles, 2010). The first
aim of the present study, therefore, was to examine the associations, among a British
sample, between 7/7 conspiracist ideation and a number of individual difference factors
previously examined by Swami et al. (2010). Specifically, we hypothesised that there
would be positive associations between 7/7 conspiracist ideation and belief in other realworld conspiracy theories, exposure to 7/7 conspiracist ideation, SDP, political cynicism,
and the Big Five personality factor of Openness to Experience, and negative associations
with attitudes towards authority and the Big Five factor of Agreeableness.
In addition, the present study also extended the work of Swami et al. (2010)
by examining the associations between 7/7 conspiracist ideation and two variables
neglected by Swami et al. (2010), namely self-esteem, and satisfaction with life. The latter
two variables were included based on Hofstadter’s (1966) argument that conspiracist
ideation is more common among the disenfranchised or disadvantaged. More specifically,
belief in conspiracy theories has been traced to self-esteem maintenance purposes
(Robins & Post, 1997; Young, 1990), wherein believers are afforded a more positive
self-image in relation to outgroups. In a similar vein, conspiracist ideation may be more
common among individuals who are dissatisfied with their lives or who believe that
malevolent actors prevent them from experiencing more fruitful lives. As such, we
expected that there would be negative associations between 7/7 conspiracist ideation
and self-esteem and satisfaction with life, respectively.
Finally, we also extended the work of Swami et al. (2010) by examining the
association between conspiracist ideation and self-assessed intelligence (SAI). This part of
the Study 1 was exploratory in nature, given that no previous work has examined associations between (either self-assessed or objective) intelligence and conspiracist ideation.
Nevertheless, there are conceptual reasons why there may be a negative association
between intelligence and conspiracist ideation. Specifically, based on the assumption that
conspiracy theories offer simplified explanations of complex events (Hofstadter, 1966;
Miller, 2002), it might be suggested conspiracy theories are more appealing to those
with lower cognitive ability. In Study 1, therefore, we examined associations between
7/7 conspiracy beliefs and SAI, which has been shown to be moderately correlated with
psychometrically measured intelligence quotient (IQ) (correlations ranging from .20 to
.50) and academic achievement (see Chamorro-Premuzic & Furnham, 2006).
Conspiracist ideation
447
Method
Participants
Participants of Study 1 were 817 (445 women, 372 men) individuals recruited from the
community in London. Participants had a mean age of 25.18 years (SD = 9.47, range
18–59 years) and the majority were of British White descent (82%; Asian = 15%, African
Caribbean = 3%). In total, 45% of participants self-reported as being atheists, 34% as
Christians, 9% as unsure of their religion, and 12% as being of some other religious
background. In terms of highest educational qualifications, 8% had obtained General
Certificates of Secondary Education (qualification generally taken by students aged 14–
16 in secondary education), 38% had obtained Advanced-Level General Certificates of
Education (part of the tertiary further education system), 42% had an undergraduate
degree, 8% had a postgraduate degree, and 4% had some other qualification.
Measures
July 7, 2005 (7/7), conspiracist ideation
To measure 7/7 conspiracist ideation, we designed a novel 12-item scale. Items for
inclusion in the scale were initially developed by the first and second authors based on
a meticulous review of the 7/7 conspiracist literature and so as to represent the range
of conspiracy theories in relation to the London bombings. Items were then carefully
worded and redundancy was eliminated by agreement between the first and second
authors. The final scale, which was rated on a 9-point Likert-type scale (1 = completely
false, 9 = completely true), is shown in Table 1. Higher scores on this scale indicate
stronger belief in 7/7 conspiracy theories. The factor structure and internal consistency
of this scale is reported in the Results section.
Belief in conspiracy theories inventory (BCTI; Swami et al., 2010)
This is a 14-item scale consisting of items describing prominent conspiracy theories.
The items were originally designed to be recognized by an international audience and
were rated on a 9-point Likert-type scale (1 = completely false, 9 = completely true).
Higher scores on this inventory indicate greater belief in a range of real-world conspiracy
theories. In their study, Swami et al. (2010) reported that this scale had a unidimensional
structure with high reliability (␣ = .86) following the exclusion of one item. In the
present study, this excluded item was replaced by a distinct item relating to conspiracist
ideation concerning the September 11, 2001, terrorist attacks, which was also previously
used by Swami et al. (2010). In the present study, Cronbach’s ␣ for the BCTI was .90.
7/7 conspiracist exposure (adapted from Swami et al., 2010)
We modified a previously used scale to measure exposure to 7/7 conspiracist ideation.
Specifically, items were adapted to refer to 7/7 conspiracy theories, rather than 9/11
conspiracy theories. The scale consists of five items describing direct (public meetings
or rallies, films or television programmes, books or articles, and websites) or vicarious
(friends or family who disbelieve official accounts of 7/7 bombings) exposure to
conspiracist ideation. The items were rated on a 9-point Likert-type scale (1 = false,
9 = true) and an overall score was computed by taking the mean of all items (higher
scores indicate greater exposure). Cronbach’s ␣ for this scale was .79, which is consistent
with previous work (Swami et al., 2010).
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Viren Swami et al.
Table 1. Items included in the 7/7 conspiracist ideation measure (Study 1) and factor loadings following
Varimax rotation
Item
9. The bombers (two recent fathers, a university graduate, and a
teacher of disabled children) do not fit the usual suicide bomber
profile, suggesting they were not the real 7/7 bombers.
10. The images of the 7/7 bombers on the days of the bombing have
been cropped and altered, suggesting that the story put out by the
police is false.
11. The initial reporting, by the police, of the bombings as simultaneous
‘power surges’ suggests that they are covering up the real cause of
the 7/7 bombings.
8. Several eyewitness accounts of the explosions on the underground
speak of explosions under the floor of the train, indicating the
bombs were planted and were not contained in backpacks.
12. British intelligence purposely failed to act on knowledge of the bomb
plot in order to ferment anti-Muslim feeling.
4. The UK government allowed the 7/7 bombings to take place so that
it would have an excuse to achieve foreign (e.g., extend wars in
Afghanistan and Iraq) and domestic (e.g., attacks on civil liberties)
goals that had been determined prior to the attacks.
7. The UK government allowed the 7/7 bombings to take place in
order to rally the country around Tony Blair’s faltering leadership.
3. The fact that the 7/7 bombers purchased return tickets and paid to
park their cars at Luton suggests that the bombings were not
planned as a suicide attack.
5. British government agencies, including the military and intelligence,
dealt incompetently with the 7/7 bombings and sought to cover up
its failures.
6. The fact that the UK government is withholding information about
the 7/7 bombings is evidence of a cover-up.
1. British intelligence had knowledge of the 7/7 bomb plot as they had
at least one of the bombers under their surveillance.
2. The 7/7 bombers were duped by British intelligence: they believed
they were carrying dummy bombs and were participating in a training
exercise testing London’s defences against backpack bombers.
Factor 1
Factor 2
.84
.12
.84
.30
.82
.39
.78
.26
.74
.28
.70
.21
.69
.34
.68
.21
.44
.83
.43
.75
.42
.70
.41
.55
Support for democratic principles (Kaase, 1971)
The SDP is a 9-item scale measuring attitudes on a number of principles common to
democratic systems, including basic democratic values. All items were rated on a 6-point
Likert-type scale (1 = complete rejection, 6 = full agreement) and following reverse
coding of several items, an overall score was computed by taking the sum of all items.
Scores ranged from 9 (least democratic) to 54 (most democratic). Cronbach’s ␣ for this
scale was .73.
Political cynicism scale (PCS; Citrin & Elkins, 1975)
This is a 13-item scale measuring concern for public interest, idealism, and political
determination, with all items on an agree-disagree format (1 = agree, 2 = disagree).
Conspiracist ideation
449
Scores are summed and higher scores indicate greater political cynicism. The internal
consistency of the scale was .82.
Attitudes to authority (AA) scale (Reicher & Emler, 1985)
Following Swami et al. (2010), we used a slightly modified, 10-item version of the AA to
measure attitudes towards authority. Specifically, participants rated, on a 5-point Likerttype scale (1 = strongly disagree, 5 = strongly agree) items relating to institutional
authority, bias by those in authority, and the absolute priority of rules, but not those
relating to the fairness of school rules. An overall score was computed by taking the mean
of all items, with higher scores reflecting more negative AA. The internal consistency of
the AA in this study was .70.
Ten-item personality inventory (TIPI; Gosling, Rentfrow, & Swann, 2003)
The TIPI is a brief scale for assessing the Big Five personality factors, which shows
adequate convergent and discriminant validity, test–retest reliability, and patterns of
external correlates (Gosling et al., 2003). Participants rated the extent to which a pair
of traits applied to them on a 7-point Likert-type scale (1 = disagree strongly, 7 = agree
strongly). Five items were reverse coded, and two items were averaged to arrive at scores
for each of the Big Five personality factors. Cronbach’s alpha coefficients were as follows:
Extraversion .54, Agreeableness .52, Conscientiousness .58, Emotional stability .50, and
Openness to Experience .51. Although these alphas are low, they were measured using
only two items and are in line with norms (for a discussion, see Gosling et al., 2003).
Rosenberg’s self-esteem scale (RSES; Rosenberg, 1965)
Self-esteem was measured using the widely used RSES, a 10-item scale that taps selfworth. Items were rated on a 4-point scale (1 = strongly disagree, 4 = strongly agree),
and an overall score was computed by taking the sum of all items, following reverse
coding of five items (higher scores indicate greater self-esteem). Scores on the RSES have
been shown to have high internal consistency and good convergent validity (Robinson
& Shaver, 1973). In the present study, Cronbach’s ␣ for the RSES was .87.
Satisfaction with life scale (SWLS; Diener, Emmons, Larsen, & Griffin, 1985)
The SWLS is a brief, 5-item measure of subjective life satisfaction that has been shown
to have good psychometric properties (e.g., Pavot & Diener, 2008). The five items are
rated on a 5-point Likert-type scale (1 = strongly disagree, 5 = strongly agree), and an
overall score is computed by taking the sum of all items (higher scores indicate greater
satisfaction with life). Cronbach’s ␣ for this scale in the present study was .83.
Self-assessed intelligence (Furnham & Gasson, 1998)
This scale presents participants with a normal IQ distribution (M = 100, SD = 15), based
on which participants were asked to estimate their overall intelligence by selecting
an actual IQ score. Scores on this scale have been associated with psychometrically
measured intelligence and academic achievement (Chamorro-Premuzic & Furnham,
2006).
450
Viren Swami et al.
Demographics
Participants provided their demographic details, consisting of sex, age, ethnicity,
religion, and highest educational qualification.
Procedure
Once ethical approval for this study was obtained from the relevant university ethics
committee, six experimenters recruited participants using a snowball sampling technique. A total of 1,000 individuals were invited to take part in the study, representing a
response rate of 81.7%. Participants were initially recruited through personal contacts of
experimenters and were asked to recruit further participants through their own networks
of contacts. Participants who agreed to take part were provided with an information
sheet about the study and provided informed consent. All participants completed
paper-and-pencil questionnaires, in which the order of the scales described above was
semi-randomized (demographics were always collected last). The questionnaires were
completed individually and anonymously, and all participants took part on a voluntary
basis. Once participants returned their completed questionnaires to their contact person,
they were provided with a debrief sheet containing further information about the study
and the experimenters’ contact details.
Results
7/7 conspiracist beliefs
To examine the factor structure of the 7/7 conspiracist ideation measure, the 12 items
were initially subjected to an exploratory factor analysis. The significance of Bartlett’s test
of sphericity, ␹ 2 (66) = 7257.07, p < .001, and the size of the Kaiser-Meyer-Olkin measure
of sampling adequacy, KMO = .92, showed that the 12 items had adequate common
variance for factor analysis (Tabachnick & Fidell, 2007). We, therefore, conducted an
exploratory factor analysis using Varimax rotation. The number of factors to be extracted
was determined by factor eigenvalues ␭ > 1.0, inspection of the scree plot (Cattell, 1966),
and an extraction criterion of .40 (Kline, 1986).
Based on the above criteria, two factors emerged with eigenvalues ␭ > 1.0 after three
iterations, with a decline between the first and second factors of the rotated solution
(␭ = 5.11 and 3.303, with 42.5% and 25.3% of the variance explained, respectively).
However, the factor loadings showed that, based on the criterion established by Kline
(1986), all items adequately loaded onto the first factor (see Table 1). Moreover, the
two factors were very highly inter-correlated (r = .84). For these reasons, we chose to
compute a single factor score for 7/7 conspiracist ideation by taking the mean of all 12
items. Cronbach’s ␣ for this measure was .93, suggesting very high internal reliability.
An independent samples t-test revealed that women were more likely to endorse 7/7
conspiracist ideation than men (women M = 3.44, SD = 1.47; men M = 3.16, SD =
1.64), t(815) = 2.62, p = .009, d = .18.
Inter-scale correlations
Descriptive statistics (M and SD) for all variables of interest are reported in Table 2.
Bivariate correlations between these variables showed that stronger belief in 7/7 conspiracy theories was significantly correlated with a stronger belief in general conspiracy
theories, greater exposure to 7/7 conspiracist ideation, greater SDP, higher political
cynicism, more negative AA, lower Agreeableness, higher Openness, lower self-esteem,
3.31
1.55
1.85
1.42
24.91
4.53
.19∗∗
.11∗
.46∗∗
3.40
1.42
.19∗∗
.50∗∗
.75∗∗
(4)
(3)
(2)
19.46
1.73
.19∗∗
.10∗
−.02
.24∗∗
(5)
2.84
.69
.33∗∗
.34∗∗
−.13∗
−.14∗
.24∗∗
(6)
4.76
1.39
−.02
.04
.03
−.02
.01
−.04
(7)
4.88
1.06
−.07∗
−.17∗
.17∗
−.06
−.33∗∗
.08∗
−.11∗
(8)
4.81
1.58
−.04
−.11∗
.23∗
−.28∗∗
−.27∗∗
.03
.40∗∗
.01
(9)
4.79
1.43
.07
−.02
−.04
−.04
−.23∗∗
.18∗
.31∗∗
.19∗∗
.03
(10)
5.15
1.17
.19∗∗
.17∗
.03
.03
.10∗
.22∗∗
.02
.03
.02
.14∗∗
(11)
31.68
5.32
−.20∗∗
−.21∗∗
.02
−.08∗
−.24∗∗
.38∗∗
.26∗∗
.32∗∗
.51∗∗
.08∗
−.16∗∗
(12)
17.24
4.04
−.21∗∗
−.13∗∗
−.02
−.25∗∗
−.24∗∗
.18∗∗
.24∗∗
.17∗∗
.28∗∗
.02
.50∗∗
−.19∗∗
(13)
116.00
12.23
−.18∗∗
−.15∗∗
−.10∗
.11∗
−.10∗
.24∗∗
−.06
−.09∗
−.05
.18∗∗
.04
−.07∗
−.18∗∗
(14)
Note. N, 817. BCTI, belief in conspiracy theories inventory; Exposure, exposure to 7/7 conspiracist ideation; SDP, support for democratic principles; PCS, Political
Cynicism Scale; AA, attitude to authority; RSES, Rosenberg’s Self-Esteem Scale; SWLS, Satisfaction with Life Scale; SAI, self-assessed intelligence. ∗ p ⬍ .05; ∗∗ p ⬍
.001.
M
SD
(1) 7/7 conspiracist
ideation
(2) BCTI
(3) Exposure
(4) SDP
(5) PCS
(6) AA
(7) Extraversion
(8) Agreeableness
(9) Conscientiousness
(10) Emotional Stability
(11) Openness
(12) RSES
(13) SWLS
(14) SAI
(1)
Table 2. Descriptive statistics and inter-scale correlations between all variables included in Study 1
Conspiracist ideation
451
452
Viren Swami et al.
Table 3. Results of the multiple regression analysis for Study 1
Variable
Belief in general conspiracy theories
Exposure to 7/7 conspiracy theories
Political cynicism
Support for democratic principles
Attitude to authority
Self-esteem
Agreeableness
Self-assessed intelligence
Satisfaction with life
Openness to experience
β
t
p
.61
.23
.13
.09
.09
−.08
−.07
−.03
−.03
.01
22.52
8.98
5.68
3.84
3.46
−3.05
−2.86
−1.41
−1.25
0.60
⬍.001
⬍.001
⬍.001
⬍.001
.001
.002
.004
.160
.213
.550
lower satisfaction with life, and lower SAI. The same pattern of correlations was also
found for belief in general conspiracy theories, suggesting that our results are robust.
Multiple regression
Although both belief in 7/7 conspiracy theories and belief in general conspiracy theories
could be considered distinct criterion variables for analytic purposes, our intention in the
present study was to predict the former. For this reason, we used belief in 7/7 conspiracy
theories as a criterion variable in a multiple linear regression (enter method). Because
three of the Big Five personality factors (Extraversion, Conscientiousness, and Emotional
Stability) did not correlate with conspiracist ideation (consistent with previous work),
we excluded these factors from our analysis. All remaining variables were included
as predictor variables, namely belief in general conspiracy theories, exposure to 7/7
conspiracy theories, SDP, political cynicism, AA, Agreeableness, Openness, satisfaction
with life, self-esteem, and SAI.
Results showed that the overall regression was significant, F(10, 816) = 134.00,
p < .001, Adj. R2 = .62. Of the variables entered into the model, the strongest
predictor was clearly belief in general conspiracy theories. In addition, exposure to 7/7
conspiracy theories, political cynicism, SDP, AA, self-esteem, and Agreeableness emerged
as significant predictors. SAI, satisfaction with life, and Openness did not emerge as
significant predictors in this analysis (see Table 3 for standardised β values, t-values, and
significance levels).
Discussion
The results of Study 1 both corroborate and extend previous work on the individual
difference correlates of belief in conspiracy theories. First, we showed that the strongest
predictor of belief in 7/7 conspiracy theories was belief in other, general conspiracy
theories. This is consistent with the suggestion that conspiracist ideation forms part of
a monological belief system (Goertzel, 1994), where one conspiratorial idea serves as
evidence for other conspiracist ideation. In the case of the present study, it might be
suggested that belief in a range of conspiracy theories provides a basis for comprehending
and accepting 7/7 conspiracy theories.
The results of Study 1 also corroborate previous work (Swami et al., 2010) in showing
significant associations between 7/7 conspiracist ideation and greater exposure to 7/7
Conspiracist ideation
453
conspiracist ideation, either directly or vicariously. Furthermore, and consistent with
previous research, our results showed significant associations between 7/7 conspiracist
ideation and greater political cynicism, more negative AA, and greater SDP. As Swami
et al. (2010) have argued, it is possible to explain this set of findings as a function of
greater anomie among conspiracy theorists. That is, 7/7 conspiracy theorists appear to
be individuals who are more disaffected with, or defiant of, the political system, but who
want greater accountability.
The results of Study 1 also showed that Agreeableness was negatively associated with
7/7 conspiracist ideation. Previously, Swami et al. (2010) explained this association as
a function of the relationship between disagreeableness and suspicion or antagonism
towards others. In our study, we also showed that Openness to Experience as positively
correlated with 7/7 conspiracist ideation, although this factor did not emerge as a
significant predictor once other factors had been accounted for. Our results also showed
that satisfaction with life and especially self-esteem was negatively associated with 7/7
conspiracist ideation. These results are consistent with the idea that conspiracist ideation
is more common among the disenfranchised or disadvantaged (Abalakina-Paap et al.,
1999; Goertzel, 1994; Hofstadter, 1966) and that conspiracy theories may be involved in
self-esteem maintenance (Robins & Post, 1997; Young, 1990).
Finally, our results also showed that there was a significant and negative correlation
between 7/7 conspiracist ideation and SAI, although the latter did not emerge as
a significant predictor in our multiple regression analysis. As we suggested above,
conspiracy theories may be more appealing to those with lower cognitive ability, given
the explanatory simplicity offered by such theories. However, an important limitation
of our design was the use of a measure of SAI. Although this measure correlates with
psychometrically measured intelligence and academic achievement (Chamorro-Premuzic
& Furnham, 2006), reported associations have tended to be moderate at best. As such,
it would be useful to explicitly examine the association between conspiracist ideation
and psychometrically measured intelligence.
STUDY 2: BELIEF IN AN ENTIRELY FICTITIOUS
CONSPIRACY THEORY
Study 2 builds on the results of Study 1 in two important ways. First, we sought to
examine whether there would be an association between belief in an entirely fictitious
conspiracy theory and belief in other, non-fictitious conspiracy theories. Specifically, we
designed a novel scale designed to tap conspiracist ideation in relation to Red Bull. Red
Bull was selected because the brand is well known in Austria (having been developed
by an Austrian) and because its parent company is one of the most successful Austrian
companies of past several decades. Both these aspects might be a good ‘breeding ground’
for a conspiracy theory, which led us to develop a novel scale tapping conspiracist
ideation that the success of Red Bull was based on a conspiracy. Evidence of an
association between belief in this entirely fictitious conspiracy theory and real-world
conspiracy theories would provide strong evidence for a monological belief system in
relation to conspiracist ideation (Goertzel, 1994).
Second, Study 2 also aimed to examine the associations between conspiracist
ideation and as yet neglected variables, namely paranormal ideation, superstitious beliefs,
psychometrically measured intelligence, and social conformity. In the first instance,
paranormal ideation refers to beliefs in phenomena outside the explanatory power of
mainstream science, such as astrology, belief in ghosts, extrasensory perception, and
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Viren Swami et al.
psychic powers (e.g., Rice, 2003). Superstitious beliefs, on the other hand, refer to
beliefs that luck or future events can be influenced by forces, rituals, or actions not
directly related to those events (e.g., Wiseman & Watt, 2004). Although previous work
has not considered possible associations between conspiracist ideation and paranormal
and superstitious beliefs, respectively, we believe there are two main reasons to expect
positive relationships.
First, conspiratorial, paranormal, and superstitious ideation may be predicated upon
a common thinking style, as each largely rejects official mechanisms of informationgeneration and expert opinion, relying instead on lay experience for legitimation (in
relation to paranormal thinking, see e.g., Irwin, 2009; Zusne & Jones, 1989). Moreover,
paranormal and superstitious beliefs have been positively associated with intuitive (rather
than analytical) thinking styles (Aarnio & Lindeman, 2005), and a similar argument may
be adequately applied to conspiracist ideation (cf. Sunstein & Vermeule, 2009).
Second, paranormal and superstitious beliefs have been shown to correlate with low
self-efficacy (e.g., Tobacyk & Shrader, 1991), external locus of control, and a tendency
to believe personal outcomes are governed by other powerful individuals, institutions,
forces, or luck (e.g., Irwin, 2009; Tobacyk, Nagot, & Miller, 1988). It might be argued,
therefore, that paranormal, superstitious, and conspiracist ideation are all predicated on
uncertainty in the face of everyday life or particular crises, which gives rise to a need for
an illusion of control. As such, we expected to find significant and positive correlations
between conspiracist ideation and paranormal and superstitious beliefs, respectively.
Study 2 also examined the association between conspiracist ideation and conformity,
which has been defined as ‘a characteristic willingness to identify with other and emulate
them, to give in to others so as to avoid negative interactions, and generally, to be a
follower rather than a leader’ (Mehrabian, 2005, p. 2). Finally, we also included measures
of self-esteem and psychometrically measured intelligence. The latter was included so as
to overcome the limitations of using a measure of SAI, as discussed in Study 1. In terms
of these factors, we expected significant and negative associations between conspiracist
ideation and self-esteem and intelligence, respectively, and a positive association with
conformity.
Method
Participants
The participants of Study 2 were 169 women and 112 men from Austria1 , all of whom
were of European White descent. Participants ranged in age from 18 to 80 years (M =
33.83, SD = 15.44). Of the total sample, 5.8% had completed primary education, 24.2%
had an apprentice diploma, 55.4% had completed secondary education, and 14.6% had
a university degree.
Materials
Entirely fictitious red bull conspiracy theory
This is a novel 12-item scale devised for the present study and consisting of items
describing completely fictititous conspiracist ideation vis-à-vis Red Bull in the Austrian
1 Twenty-three
percent of participants were from southern parts of Germany, which are culturally very similar to Austria.
Conspiracist ideation
455
Table 4. Items included in the entirely fictitious conspiracy theory measure (Study 2) and factor
loadings following Varimax rotation
Initial factor loadings
Item
7. Regular consumption of Red Bull raises dopamine levels,
which causes damage in the long term.
9. In the beginning, Red Bull was illegal for minors in Austria,
which raises questions as to its subsequent legalisation.
10. Commercials in sports give the impression that Red Bull is
healthy.
8. In 2001, the 23-year-old man Klaus Weber died of cerebral
haemorrhage, caused by overly high consumption of Red
Bull.
2. Red Bull contains illegal substances that raise the desire for
the product
4. If a can of Red Bull is heated up to 40◦ C, it releases
health-threatening substances.
5. Subliminal messages in Red Bull television commercials
make consumers believe that Red Bull improves one’s
health.
11. The extract ‘testiculus taurus’ found in Red Bull has
unknown side effects.
6. The slogan ‘Red Bull gives you wings’ is used because in
animal experiments, rats grew rudiment wings.
1. The recipe of Red Bull, originally used as doping for
soldiers, was bought from an American officer.
3. The official inventor of Red Bull, Dietrich Mateschitz, pays
10 million Euros each year to keep food controllers quiet.
12. The ban of Red Bull Cola in Germany was due to not
paying enough bribe money to German politicians.
Factor 1
Factor 2
Final factor
loadings
.78
.16
.75
.75
.15
.71
.75
−.07
.55
.70
.27
.69
.69
.38
.75
.68
.39
.69
.65
.36
.67
.57
.40
.68
−.15
.79
-
.41
.65
.61
.50
.58
.73
.45
.57
.69
context. The 12 items were initially developed by research assistants who were familiar
with the nature of the study. Items were then critiqued and revised to eliminate
redundancy and ensure ease of understanding. All items represent entirely fictititous
aspects of a Red Bull conspiracy theory. The final list of items is reported in Table 4.
Participants rated the extent to which they agreed that each of the final statements was
true or false on a 9-point scale (1 = completely false, 9 = completely true), with higher
scores indicating greater belief in this fictitious conspiracy theory. The factor structure
and reliability of this scale are reported in the Results section.
Belief in conspiracy theories inventory (Swami et al., 2010)
This scale was identical to that used in Study 1. Cronbach’s ␣ for this scale in Study 2
was .87.
Australian sheep-goat scale (ASGS; Lange & Thalbourne, 2002; German translation: Voracek, 2009)
The ASGS is an 18-item measure of paranormal beliefs and experiences, where ‘sheeps’
refer to those who believe in the possibility of paranormal phenomena and ‘goats’
refer to disbelievers. Following Voracek (2009) and for consistency, the visual analogue
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Viren Swami et al.
system format of the original ASGS was modified to a 6-point scale (0 = totally disagree,
5 = totally agree). Although the scale may contain three different subscales (relating
to extrasensory perception, psychokinesis, and belief in the afterlife), subscale scores
are highly correlated and most researchers (including those who developed the scale)
derive an overall mean representing belief in paranormal phenomena (e.g., Thalbourne,
1995a, 1995b; Thalbourne, Dunbar, & Delin, 1995; Voracek, 2009). Cronbach’s ␣ for
the ASGS in the present study was .94, which is identical to a previous study using this
version of the scale (Voracek, 2009).
Superstitious beliefs (Wiseman & Watt, 2004; German translation: Voracek, 2009)
Superstitious beliefs were measured using a 6-item scale developed by Wiseman and
Watt (2004), which measures agreement with both negative (e.g., walking under a
ladder) and positive (e.g., touching wood) superstitions (0 = totally disagree, 5 = totally
agree). Although it is possible to derive scores for negative and positive superstitious
beliefs separately, previous work using the German version of this scale has shown
that it best considered unidimensional (Voracek, 2009). In the present study, therefore,
we computed an overall superstitious beliefs score by taking the mean of all 6 items.
Cronbach’s ␣ for this scale in the present study was .86, which was similar to that
reported in a previous study (Voracek, 2009).
Wortschatztest (German Vocabulary Test, WST; Schmidt & Metzler, 1992)
This is multiple-choice vocabulary test that provides a measure of crystallized (general)
intelligence. The measure comprises 42 items, each of which consists of one target
word plus five pseudo-word distractors. An overall score is computed by taking the
sum of correctly answered items. The WST is frequently used for estimating cognitive
functioning among German-speaking samples and has been shown to be correlated with
other measures of cognitive ability (e.g., Merten, 2005).
Conformity scale (CS; Mehrabian, 2005)
This 11-item scale measures conformity on a 9-point scale (−4 = very strong disagreement, + 4 = very strong agreement). A total score is computed by summing
participants’ responses to seven positively worded items and by subtracting this value
from the sum of their responses to four negatively worded items. Because of this scoring
technique, it is not possible to calculate Cronbach’s ␣ for this scale, although previous
work has shown that the scale has good test–retest reliability and construct validity
(Mehrabian & Stefl, 1995).
Rosenberg self-esteem scale (RSES; Rosenberg, 1965)
This scale was identical to the one used in Study 1, with the exception of coding (0 =
strongly disagree, 3 = strongly agree). In the present study, Cronbach’s ␣ for this scale
was .77.
Demographics
Participants provided their demographic details consisting of sex, age, and highest
educational qualification.
Conspiracist ideation
457
Procedure
Unless stated above, three researchers (the third, fourth, and final authors) developed
German translations of the scales listed above using the parallel blind technique
(Behling & Law, 2000). This involved each researcher translating the questionnaire
by himself. The translations were then compared against each other and disputed
phrases or translations were settled by agreement between the researchers. Following
questionnaire preparation, a multitude of data collectors directly recruited participants
through their personal contacts using a snowball-sampling technique. As in Study 1,
several experimenters each recruited participants through their personal contacts, who
in turn were requested to recruit further participants through their own networks.
All participants took part on a voluntary basis and were not remunerated for their
participation. Participation was entirely anonymous and confidential, and all participants
were debriefed once they had returned their questionnaire to the researchers.
Results
Entirely fictitious conspiracy theory
To examine the factor structure of the Red Bull conspiracy theories measure, the 12
items were initially subjected to an exploratory factor analysis. The significance of
Bartlett’s test of sphericity, ␹ 2 (66) = 1264.44, p < .001, and the size of the KaiserMeyer-Olkin measure of sampling adequacy, KMO = .90, showed that the 12 items
had adequate common variance for factor analysis (Tabachnick & Fidell, 2007). We,
therefore, conducted an exploratory factor analysis using Varimax rotation, and using
the same criteria as delineated in Study 1.
Our results showed that there were two factors with eigenvalues ␭ > 1.0 after
three iterations, with a steep decline between the first and second factors of the rotated
solution (␭ = 5.27 and 1.17, with 43.9% and 9.7% of the variance explained, respectively).
The factor loadings showed that only one item did not load onto the first factor (item 6)
(see Table 4). We, therefore, dropped this item and repeated the factor analysis. Results
showed that there was a unidimensional solution with the 11 items, with ␭ = 5.17
and 47.0% of the variance explained (see Table 4 for factor loadings). Based on these
results, we computed an overall score to represent agreement with the entirely fictitious
conspiracy theory by taking the mean of the 11 retained items. Cronbach’s ␣ for this
measure was .89. An independent samples t-test showed that women were more likely
to endorse this entirely fictitious conspiracy theory than men (women M = 3.88, SD =
1.47; men M = 3.48, SD = 1.43), t(276) = 2.26, p = .025, d = .27.
Inter-scale correlations
Descriptive statistics for all variables (belief in the entirely fictitious conspiracy theory,
belief in general conspiracy theories, paranormal beliefs, superstitious beliefs, crystallised
intelligence, conformity, and self-esteem) are reported in Table 5. Bivariate correlations
between these variables showed that stronger belief in the entirely fictitious conspiracy
theory was significantly correlated with a stronger belief in general conspiracy theories,
greater paranormal beliefs, greater superstitious beliefs, and lower self-esteem, but not
with any of the other variables. In addition, a stronger belief in general conspiracy
theories was significantly correlated with greater paranormal beliefs, lower self-esteem,
and lower crystallised intelligence.
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Viren Swami et al.
Table 5. Inter-scale bivariate correlations between all variables included in Study 2
(1)
(1) Red Bull conspiracy theories
(2) BCTI
(3) ASGS (paranormal beliefs)
(4) Superstitious beliefs
(5) WST (crystallised intelligence)
(6) Conformity Scale
(7) Self-esteem
M
SD
3.72
1.46
(2)
(3)
(4)
(5)
(6)
(7)
.50∗∗
.31∗∗
.37∗∗
.15∗
.11
.26∗∗
−.03
−.19∗∗
−.23∗∗
−.18∗∗
.02
.06
.02
.16∗∗
−.09
−.12∗
−.12∗
.05
−.07
.06
−.32∗∗
31.05
5.52
6.82
11.37
24.52
4.25
3.90
1.43
1.47
1.10
0.89
1.05
Note. N, 281; ∗ p ⬍ .05; ∗∗ p ⬍ .001. BCTI, belief in conspiracy theories inventory; WST, Wortschatztest;
ASGS, Australian Sheep-Goat Scale.
Table 6. Results of the multiple regression analysis for Study 2
Variable
Belief in general conspiracy theories
Paranormal beliefs
Crystallised intelligence
Self-esteem
Superstitious beliefs
Conformity
β
t
p
.45
.15
−.11
−.09
.08
−.05
8.08
2.57
−2.02
−1.61
1.51
−0.82
⬍.001
.011
.044
.109
.131
.415
Multiple regressions
As in Study 1, we computed a multiple linear regression (enter method) with belief in
the entirely fictitious conspiracy theory as the criterion variable and belief in general
conspiracy theories, paranormal beliefs, superstitious beliefs, crystallised intelligence,
social conformity, and self-esteem as the predictor variables. Results showed that the
overall regression was significant, F(6, 277) = 18.59, p < .001, Adj. R2 = .28. Of the
variables entered into the model, the only significant predictors were belief in general
conspiracy theories, paranormal beliefs, and crystallised intelligence (see Table 6 for β
values, t-values, and significance levels).
Discussion
The results of this study showed that the strongest predictor of belief in the entirely
fictitious conspiracy theory was belief in other real-world conspiracy theories. This
provides additional support for Goertzel’s (1994) argument that a person who believes in
one conspiracy theory is more likely to believe in others, including entirely fictitious ones
perceived as real. In addition, however, the results of Study 2 showed that belief in the
Red Bull conspiracy theory was predicted by paranormal beliefs. As we suggested above,
this association may be predicated on the fact that both conspiracist and paranormal
ideation are underpinned by similar thinking styles. Specifically, both these beliefs
Conspiracist ideation
459
may rely on intuitive cognisance or lay experience for legitimation, discounting official
sources of information when evaluating specific claims.
Alternatively, it might be argued that, for some individuals, the uncertainty caused by
crises results in a need to reassert control, making such occurrences comprehensible,
and potentially controllable (Baigent, Leigh, & Lincoln, 1987). In this sense, it might
be argued that conspiracist ideation provides an illusion of control, which is consistent
with recent experimental work suggesting that participants who lack control are more
likely to believe in conspiracy theories (Sullivan, Landau, & Rothschild, 2010; Whitson &
Galinsky, 2008). The association between conspiracist ideation and paranormal beliefs
thus becomes explicable, given that the latter has been associated with low self-efficacy
(e.g., Tobacyk & Shrader, 1991), external locus of control, and a tendency to believe
personal outcomes are governed by other powerful individuals, institutions, forces, or
luck (e.g., Irwin, 2009; Tobacyk et al., 1988).
The results of this study also extend the work in Study 1 by showing that Red Bull
conspiracist beliefs were significantly associated with lower crystallized intelligence.
We suggest that, to the extent that conspiracy theories offer simplified explanations of
complex phenomena (Hofstadter, 1966; Miller, 2002), such beliefs may be more readily
accepted by individuals with lower cognitive ability. However, it should also be noted
that the association between conspiracist ideation and cognitive ability was relatively
weak. Even so, this represents a novel area of research, and future work could extend the
present results by more closely examining the association between conspiracist ideation
and cognitive ability.
Our results suggest that there was no significant association between conspiracist
ideation and social conformity. This is an interesting finding, given both the results
of Study 2 and previous work (Swami et al., 2010) showing that greater exposure
to conspiracist ideation (including vicarious exposure) is associated with stronger
conspiracist beliefs. It may be the case that conspiracist ideation begins as an individual
decision-making process, although vicarious experience intensifies the move towards
such ideation. Finally, the results of Study 2 also showed that superstitious beliefs were
significantly correlated with conspiracist ideation, although this factor did not reach
significance in our regression analyses.
GENERAL DISCUSSION
Taken together, the results of the present studies suggest that conspiracist ideation
may initially begin as an individual process, in which a person tries to makes sense
of some event perceived as threatening or calamitous. In such a scenario, a tendency
towards conspiracist ideation may tend to be more prevalent among individuals who are
politically cynical, show stronger SDP, have lower self-esteem, are more disagreeable,
and possibly have lower crystallised intelligence. Systemic factors, such as discrepancies
or ambiguities in mainstream explanations for an event, may also play a role in initially
shaping conspiracist ideation (Hardin, 2002; Sunstein & Vermeule, 2009).
Once this process has been initiated, a confirmation bias and avoidance of cognitive
dissonance may further the drive towards conspiracist ideation (Douglas & Sutton,
2008). However, our results also suggest that the strongest predictor of whether or
not an individual will ultimately accept a conspiracy theory is the presence of earlier
conspiracist ideation. This is entirely consistent with Goertzel’s (1994) suggestion that
conspiracy theories form part of a monological belief system, where conspiracist ideation
460
Viren Swami et al.
increases the chances that an individual will accept evidence of novel conspiracy
theories. Such a system may allow individuals to easily comprehend new phenomena
within existing belief systems, but communal reinforcement may also play a role in
embedding conspiracy theories within particular social groups.
Interestingly, the results of the two studies here suggest this effect may also extend to
belief in entirely fictitious conspiracy theories. That is, believing in real-world conspiracy
theories appears to make it more likely that an individual will also be more accepting
of fictitious conspiracy theories. In addition, the two studies reported here showed that
women were more likely to endorse conspiracist ideation than men. This result may be
explicable in terms of the specific findings of Study 2. Previous work has shown that
women are more likely to hold paranormal beliefs than men (e.g., Rice, 2003; Vyse,
1997), which has been explained as a function of women thinking less analytically
and more intuitively than men (Aarnio & Lindeman, 2005). Clearly, sex differences in
conspiracist ideation are an area that requires further research.
The importance of the two studies reported here is that they identify a constellation
of individual difference traits that are associated with conspiracist ideation. This, in
turn, may assist policy makers identify instances where harmful conspiracy theories may
develop. For example, some scholars have noted the positive roles played by conspiracy
theories (e.g., offering individuals an opportunity to contest the credibility of sociopolitical actors), but also highlight the fact that conspiracy theories are often limited
because their critiques of existing power structures are often highly simplistic (Fenster,
1999), succumbing to racist or exclusionary politics. In such cases, a focus on individual
differences may be useful in assisting to stem harmful conspiracy theories at the root.
The main limitations of the two studies reported here are the correlational design,
which means that causal relations cannot be clarified and the non-randomized samples,
which means that results can only be generalized with caution. In addition, there are
further individual difference traits that could legitimately be examined in future work,
including locus of control, just world beliefs, and subjective happiness (Swami & Coles,
2010). Future research would also do well to more explicitly understand the context in
which conspiracy theories arise; a task that may be more suited to qualitative research
methods. Doing so will undoubtedly provide a more rounded picture of the functions
that conspiracy theories serve as well as their effects (Waters, 1997).
Acknowledgements
We are grateful to Lisa Bower and Elisabeth Farr for assistance with data collection.
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Received 18 June 2010; revised version received 24 September 2010
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